Taking Uncertainty Seriously: Bayesian Marginal Structural Models for Causal Inference
A. Jordan Nafa and Andrew Heiss
University of North Texas and Georgia State University
September 16th, 2022
Introduction
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In many fields of political science baseline random assignment and experimental design are impossible
For many research questions, observational data is our only option but this makes causal inference difficult
Assumption of strict exogeneity is usually unrealistic and valid instruments are a rare beast (Liu, Wang, and Xu 2022; Mogstad and Torgovitsky 2018; Swamy, Tavlas, and Hall 2015)
Need to rely on weaker assumptions for causal identification (Acharya, Blackwell, and Sen 2016; Blackwell and Glynn 2018; Forastiere, Mattei, and Ding 2018)
How can we estimate causal effects and describe their uncertainty using observational data?
Causal Inference in Political Science
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Cross-sectional time series data and causal inference
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Causal inference literature in political science focuses largely on frequentist econometrics (i.e., Imai and Kim 2019, 2020)
- Still relies on strict exogeneity assumption
Recent works drawing on approaches developed in biostatistics outline a framework for estimating causal effects under the relatively weaker assumption of sequential ignorability (Acharya, Blackwell, and Sen 2016; Blackwell and Glynn 2018)
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Our goal in this paper is to extend the approach to causal inference under selection on observables introduced by Blackwell and Glynn (2018) to a Bayesian framework
Marginal Structural Models
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Marginal structural models (MSMs) are a multi-stage approach to estimating causal effects where baseline random assignment is not possible (Robins 1997; Robins, Hernán, and Brumback 2000)
Relies on inverse probability of treatment weighting to achieve covariate balance by constructing pseudo-populations (Cole and Hernán 2008; Imai and Ratkovic 2015; Saarela et al. 2015)
Adjusting for biasing paths in the propensity model allows for identification of causal effects that are impossible to estimate in a single model due to post-treatment bias
Possible to estimate lagged effects and “treatment histories” in cross-sectional time series data under complex temporal dependence (Blackwell and Glynn 2018)
Why Bayesian Estimation?
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Frequentist uncertainty estimates are based on assumptions about sampling distributions
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Yet, in many areas of political science our data comprise an apparent population rather than a random sample from a larger population of interest (Berk, Western, and Weiss 1995; Gill 2001)
- For example, all sovereign countries between 1945 and 2020 or all states in U.S. over some time period
It doesn’t make sense to think in terms of random samples from a population if your observed data is the population (Gill and Heuberger 2020; Western and Jackman 1994)
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Why Bayesian Estimation?
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A Bayesian framework provides a straightforward approach to accounting for and propagating uncertainty in the specification of the propensity model
Bayesian Model Averaging (BMA) and cross-validation based stacking approaches allow us to avoid choosing a single specification for the propensity model (Kaplan and Chen 2014; Corwin Matthew Zigler and Dominici 2014)
Acknowledges that we are virtually always uncertain about what the true set of confounders is
May help reduce the degree to which our results depend on the propensity model being correctly specified (Hahn, Murray, and Carvalho 2020)
Why Bayesian Estimation?
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Inverse probability of treatment weights are an estimated quantity with associated uncertainty (Liao and Zigler 2020)
Most applications of IPTW methods in political science assume the weights are deterministic (i.e., Blackwell and Glynn 2018; Kurtz and Lauretig 2021; Ladam, Harden, and Windett 2018)
Need to propagate uncertainty in the design stage weights to outcome stage (Liao and Zigler 2020)
Highlights a problem with fully Bayesian estimation of MSMs which requires the models be estimated jointly (Robins, Hernán, and Wasserman 2015; Corwin M. Zigler et al. 2013)
A Bayesian Pseudo-Likelihood Approach
\[ \definecolor{treat}{RGB}{27,208,213} \definecolor{outcome}{RGB}{98,252,107} \definecolor{baseconf}{RGB}{244,199,58} \definecolor{covariates}{RGB}{178,26,1} \definecolor{index}{RGB}{37,236,167} \definecolor{timeid}{RGB}{244,101,22} \definecolor{mu}{RGB}{71,119,239} \definecolor{sigma}{RGB}{219,58,7} \newcommand{normalcolor}{\color{white}} \newcommand{treat}[1]{\color{treat} #1 \normalcolor} \newcommand{resp}[1]{\color{outcome} #1 \normalcolor} \newcommand{conf}[1]{\color{baseconf} #1 \normalcolor} \newcommand{covar}[1]{\color{covariates} #1 \normalcolor} \newcommand{obs}[1]{\color{index} #1 \normalcolor} \newcommand{tim}[1]{\color{timeid} #1 \normalcolor} \newcommand{mean}[1]{\color{mu} #1 \normalcolor} \newcommand{vari}[1]{\color{sigma} #1 \normalcolor} \]
Bayesian Design Stage Estimation
For some binary treatment \(\treat{X}_{\obs{i}\tim{t}}\), the posterior expectation of the stabilized inverse probability of treatment weights for each unit \(\obs{i}\) at time \(\tim{t}\) is
\[ \text{IPW}_{\obs{i}\tim{t}} = \prod^{\tim{t}}_{\tim{t} = \tim{1}} \frac{\int\Pr[\treat{X}_{\obs{i}\tim{t}}~ | ~\treat{X}_{\obs{i}\tim{t-1}},~ \conf{C}_{\obs{i}}]\pi(\theta)d\theta}{\int\Pr[\treat{X}_{\obs{i}\tim{t}}~ |~\covar{Z}_{\obs{i}\tim{t}}, ~ \treat{X}_{\obs{i}\tim{t-1}},~ \resp{Y}_{\obs{i}\tim{t-1}},~ \conf{C}_{\obs{i}}]\pi(\theta)d\theta} \]
\(\treat{X}_{\obs{i}\tim{t-1}}\) and \(\resp{Y}_{\obs{i}\tim{t-1}}\) denote the treatment status and outcome for the \(\obs{i^{th}}\) unit in the previous period respectively
\(\conf{C}_{\obs{i}}\) is a set of time-invariant baseline covariates
\(\covar{Z}_{\obs{i}\tim{t}}\) is a set of time-varying covariates that satisfies sequential ignorability
Although we focus mainly on the average treatment effect at times \(\tim{t}\) and \(\tim{t-1}\), it is possible to estimate longer lags and other estimands as well.
Bayesian Design Stage Estimation
It is also possible to extend IPTW to cases in which \(\treat{X}_{\obs{i}\tim{t}}\) is continuous, in which case the stabilized weights are
\[\text{IPW}_{\obs{i}\tim{t}} = \prod^{\tim{t}}_{\tim{t} = \tim{1}} \frac{f_{\treat{X}_{\obs{i}\tim{t}} | \treat{X}_{\obs{i}\tim{t-1}},\conf{C}_{\obs{i}}}[(\treat{X}_{\obs{i}\tim{t}}~ | ~\treat{X}_{\obs{i}\tim{t-1}},~ \conf{C}_{\obs{i}}); ~\mean{\mu}, ~\vari{\sigma^{2}}]}{f_{\treat{X}_{\obs{i}\tim{t}} |\covar{Z}_{\obs{i}\tim{t}}, \treat{X}_{\obs{i}\tim{t-1}}, \resp{Y}_{\obs{i}\tim{t-1}}, \conf{C}_{\obs{i}}}[(\treat{X}_{\obs{i}\tim{t}}~ |~\covar{Z}_{\obs{i}\tim{t}}, ~ \treat{X}_{\obs{i}\tim{t-1}},~ \resp{Y}_{\obs{i}\tim{t-1}},~ \conf{C}_{\obs{i}}); ~\mean{\mu}, ~\vari{\sigma^{2}}]} \]
Each of the parameters \(\treat{X}\), \(\resp{Y}\), \(\covar{Z}\), and \(\conf{C}\) in the numerator and denominator are the same as in the binary version
The \(f_{\dots}(\cdot)\) expressions represent a probability density function with mean \(\mean{\mu}\) and variance \(\vari{\sigma^{2}}\)
We’ll focus mainly on binary treatment regimes, though this particular method tends to behave better for a continuous \(\treat{X}\) in some cases
The Bayesian Pseudo-Likelihood
To propagate uncertainty in the distribution of weights from the design stage while avoiding the problem of feedback inherent in joint estimation, we develop a Bayesian Pseudo-Likelihood estimator (Savitsky and Toth 2016; Williams and Savitsky 2020a, 2020b)
\[\begin{align} \hat{\pi}( \theta~|~y, \tilde{w}) ~\propto~ \left [\prod_{i = 1}^{n} \Pr(y_{i} ~|~ \theta)^{\tilde{w_{i}}}\right ]\pi(\theta) \end{align}\]\(\tilde{w_{i}}\) is the realized IPT weight for the \(i^{th}\) observation
\(\prod_{i = 1}^{n} \Pr(y_{i} ~|~ \theta)^{\tilde{w_{i}}}\) is the pseudo-likelihood and \(\pi\) denotes the prior probability for a parameter \(\theta\)
\(\hat{\pi}( \theta~|~y, \tilde{w})\) represents the Bayesian pseudo-posterior for \(\theta\)
Parameterization of the Weights
We decompose the matrix of weights from the design stage into a location component \(\lambda\) and a scale component \(\delta\)
The weight for each observation is sampled as \[\tilde{w}_{\obs{i}\tim{t}} \sim \lambda_{\obs{i}\tim{t}} + \delta_{\obs{i}\tim{t}} \cdot \pi(\delta_{\obs{i}\tim{t}})\] where \(\pi(\delta_{\obs{i}\tim{t}})\) is a regularizing prior on the scale of the weights such as an exponential distribution with rate \(\lambda > 3.5\) or Beta distribution with shape parameters \(\alpha = 2\) and \(\beta \ge 2\)
Provides computational stability and shuts down extreme values when the IPT weights have high variance
Straightforward extensions for nested data structures via double-weighted estimation (Savitsky and Williams 2021)
Simulation Study Design
Simulation Study Overview
To assess parameter recovery and bias, we adapt the original simulation design from Blackwell and Glynn (2018)
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We simulate 2000 data sets of varying dimensions, manipulating the path \(\treat{X}_{\obs{i}\tim{t-1}} \longrightarrow \covar{Z}_{\obs{i}\tim{t}}\)
Periods \(\in \{20, 50\}\)
Groups \(\in \{25, 45, 65, 85, 100\}\)
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Objectives
Identify both \(\treat{X}_{\obs{i}\tim{t}} \longrightarrow \resp{Y}_{\obs{i}\tim{t}}\) and \(\treat{X}_{\obs{i}\tim{t-1}} \longrightarrow \resp{Y}_{\obs{i}\tim{t}}\)
Compare our Bayesian Pseduo-Likelihood approach against the more common auto-regressive distributed lag (ARDL) specification
DAG for the Simulated Data
ARDL Model Specification
\[\begin{align} \resp{y}_{\obs{i}\tim{t}} &\sim \textit{Normal}(\mu_{\obs{i}\tim{t}}, \epsilon^{2})\\ &\mu_{\obs{i}\tim{t}} = \alpha + \beta_{1}\treat{X}_{\obs{i}\tim{t}} + \beta_{2}\treat{X}_{\obs{i}\tim{t-1}} + \beta_{3}\resp{Y}_{\obs{i}\tim{t-1}} + \beta_{4}\resp{Y}_{\obs{i}\tim{t-2}} +\\ & \quad \beta_{5}\covar{Z}_{\obs{i}\tim{t}} + \beta_{6}\covar{Z}_{\obs{i}\tim{t-1}} + \epsilon\\ \text{with priors}\\ \alpha &\sim \textit{Normal}(\bar{y}, ~ 2 \cdot \sigma_{y}) \quad\quad\quad \beta_{k} \sim \textit{Normal}\left(0, ~ 1.5 \cdot \frac{\sigma_{y}}{\sigma_{x}}\right)\\ \epsilon &\sim \textit{Exponential}\left(\frac{1}{\sigma_{y}}\right) &\\ \end{align}\]MSM Design Stage Specification
As illustrated in the equation for the stabilized weights, we specify two separate models for the numerator and denominator with weakly informative independent normal priors on \(\alpha\) and \(\beta\)
\[\begin{align} \Pr(\treat{X}_{\obs{i}\tim{t}} = 1 ~|~ \theta_{\obs{i}\tim{t}}) &\sim \textit{Bernoulli}(\theta_{\obs{i}\tim{t}})\\ &\theta_{\obs{i}\tim{t}} = \text{logit}^{-1}(\alpha + X_{n}\beta_{k})\\ \text{with priors}\\ \alpha &\sim \textit{Normal}(0, ~2) \quad \quad \beta_{k} \sim \textit{Normal}(0,~ 1)\\ \end{align}\]For the numerator model, the matrix \(X_{n}\) is simply \(\treat{X}_{\obs{i}\tim{t-1}}\)
For the denominator model, \(X_{n} = \{\covar{Z}_{\obs{i}\tim{t}}, ~ \treat{X}_{\obs{i}\tim{t-1}},~ \resp{Y}_{\obs{i}\tim{t-1}}\}\)
MSM Outcome Model Specification
\[\begin{align} \resp{y}_{\obs{i}\tim{t}} &\sim \textit{Normal}(\mu_{\obs{i}\tim{t}}, \epsilon^{2})^{\tilde{w}_{\obs{i}\tim{t}}}\\ &\mu_{\obs{i}\tim{t}} = \alpha + \beta_{1}\treat{X}_{\obs{i}\tim{t}} + \beta_{2}\treat{X}_{\obs{i}\tim{t-1}} + \epsilon & \\ \text{where}\\ \tilde{w}_{\obs{i}\tim{t}} &\sim \lambda_{\obs{i}\tim{t}} + \delta_{\obs{i}\tim{t}} \cdot \pi{(\delta)}\\ \text{with priors}\\ \alpha &\sim \textit{Normal}(\bar{y}, ~ 2 \cdot \sigma_{y}) \quad \quad \beta_{k} \sim \textit{Normal}\left(0, ~ 1.5 \cdot \frac{\sigma_{y}}{\sigma_{x}}\right)\\ \epsilon &\sim \textit{Exponential}\left(\frac{1}{\sigma_{y}}\right) \quad \quad \delta_{\obs{i}\tim{t}} \sim \textit{Beta}(2, ~ 5)\\ \end{align}\]Simulation Results
Simulation Results
Simulation Results
Simulation Results
Simulation Results
Simulation Results
Conclusions
Conclusion
Overall, our proposed procedure performs well in terms of parameter recovery under fairly general conditions
Going forward, we need to apply this to some real world political science examples
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Planned R package implementing our procedure by building on the brms package as a back end
- Makes it super easy for anyone who knows standard R model syntax to use